N = 676). Results indicated the scale was unidimensional and demonstrated excellent psychometric properties. The impact of memory characteristics on symptoms of Prolonged Grief Disorder (PGD) according to both ICD-11 and DSM-5-TR criteria were investigated using cross-lagged structural equation modelling in a three-wave longitudinal sample (N = 275) at baseline and 6 and 12 months later. Results indicated that loss-related memory characteristics predicted future symptoms of PGD after controlling for autoregressions, and concurrent associations between symptoms and memory characteristics. Cross-lagged associations between memory characteristics and symptoms were significant in the first 6 months of follow-up. After that, memory characteristics predicted future symptoms, but not the other way round. Theoretical and clinical utility of the scale and its features are discussed.Difficulties with loss-related memories are hypothesised to be an important feature of severe and enduring grief reactions according to clinical and theoretical models. However, to date, there are no self-report instruments that capture the different aspects of memory relevant to grieving and adaptation after bereavement over time. The Oxford Grief–Memory characteristics scale (OG-M) was developed using interviews with bereaved individuals and was subject to exploratory and confirmatory factor analyses in a community sample ( Following a bereavement, severe and enduring grief is thought to persist in approximately 7–14% of grievers . In receTheoretical models of prolonged grief hypothesise that symptoms of grief result from a failure to integrate information about the reality of the loss into the autobiographical memory base –8. ShearPosttraumatic Stress Disorder (PTSD), like prolonged grief, can also be conceptualised as a disorder that arises from a failure of memory integration of an extremely upsetting event , 11. ForPrevious investigations of the characteristics of memory associated with severe and enduring grief have focused on the content of intrusive imagery  and over“When I try to remember good times we have shared, memories of the loss pop up”) giving rise to predominantly negative emotions . Distressing memories were reported as easily triggered  and activated in response to specific cues such as seeing others with their partners or families. Prior memories of the past and experiences with the deceased highlighted the incongruence of self-image since the loss  and memories without the deceased both positive and neutral were less accessible . Memories of the loss were also connected to the emotions felt at the time of the loss  and had a sense of “nowness” . Memories also involved a sense of visceral changes in the bereaved individual’s body in the form of pain , or physical deficit .The content of the interviews informed a series of items aimed to measure multiple features of memory associated with symptoms of PGD. These include descriptions of loss-related memories as intrusive and distressing in people with PGD , as wellTo date, there is no self-report scale that captures the memory characteristics that may be associated with symptoms of prolonged grief. Such a tool could aid clinicians in their decision-making by highlighting the characteristics of memory associated with prolonged grief that need to be targeted in treatment. The scale could also perform as a process measure allowing investigation of potential mediation effects to better understand clinical outcomes. Finally, the scale could inform adaptations to treatment for individual patients. For example, if a patient endorsed high triggering of intrusive memories by many cues in everyday life, this would indicate clinical intervention focused on stimulus discrimination prior to loss memory work. Previous research has highlighted the importance of exposure to distressing memories in treating prolonged grief  but to oTo address this gap, we present a new loss-specific measure, The Oxford Grief Memory Characteristics Scale (OG-M). Developed from interviews described in Smith, Rankin and Ehlers , this inThree separate samples of bereaved individuals were recruited via social media advertisements, bereavement charity mailing lists, and the Google content network in the United Kingdom. Demographics and loss characteristics for each sample are described in A third three-wave longitudinal sample assessed in the first 6 months after bereavement and again 6 and 12 months later consisted of 275 adults and was used to assess the effect of the scale over time.Questionnaire data were collected online and followed an ethical framework for internet research . All parI am reminded of the loss for no apparent reason”), qualities of memory , the poor availability of positive memories , and the physical impact of loss-related memories . Four further items, taken from the Intrusions Questionnaire   was usedThe 10 symptom DSM-5-TR criteria can be fully represented with these items . For theThe PCL-5 is a self-report instrument assessing distress associated with the 20 symptoms of PTSD in DSM-5 over the past month. Items were rated on a five-point scale, from . Internal consistency was excellent in all samples .The PHQ-9 is a self-report measure that mirrors the nine major depressive symptoms for major depressive disorder according to the Diagnostic and Statistical Manual, 4th Edition, Text Revision [DSM IV-TR ]. Each i“I feel compelled to touch things that they touched ”], activities ”] and experiences  aimed at restoring or maintaining proximity to the deceased person. Composite reliability was good in the cross-sectional sample (ω = 0.87).This 23-item questionnaire asks participants on a 5-point scale  to indicate how often they used particular strategies to cope with their loss. Items pertain to four content factors: Avoidance, Proximity seeking, Grief rumination, and Injustice rumination. The 7-item proximity seeking subscale measures the extent to which bereaved individuals engage in behaviours [e.g., 2 goodness-of-fit test where the χ2:df ratio is smaller than 3:1 was considered acceptable. A comparative fit index (CFI) or Tucker Lewis Index (TLI) of 0.90 or higher or 0.95 or higher was considered acceptable and good respectively. For root mean square error of approximation (RMSEA) 0.08 or lower was considered acceptable and 0.06 or lower was considered good , where λi are the factor loadings and δii, the error variances) for WLSMV on the total scale in the CFA  the CFA . Criteri the CFA . A score the CFA .Next, we aimed to create a clinician-version consisting of a shortened list of items from the original memory characteristics scale that balanced predictive validity with ease of use. The OG-M items were subject to Area Under the Curve (AUC) analysis of Receiver Operating Characteristics curves (ROC) . Each itFinally, in order to determine the predictive role of memory characteristics on grief severity we employed a second-order autoregressive cross-lagged panel model in Mplus Version 8 . We pres2p > 0.05, CFI > 0.90, TLI > 0.90, RMSEA < 0.01 (A full information maximum likelihood (FIML) approach implemented in Mplus was used to estimate missing data to minimise the bias associated with attrition. A majority of participants completed the OG-M at all three time points (66.2%) with a high proportion answering at least two (87.6%); no participants were excluded. Covariance coverage, which measures the impact of missing data, ranged from 0.66 to 0.93 for each pair of variables, well above the minimum threshold of 0.10 for model convergence . The folA < 0.01 –36.2 = 636.55 on df = 324, χ2:df = 1.97). The two-factor solution fit statistics indicated a slightly worse fit to the data , χ2 = 664.72 on df = 298, χ2:df = 2.23) and the three-factor solution, while demonstrating the best fit , χ2 = 480.51 on df = 273, χ2:df = 1.76), was not conceptually interpretable. No items loaded strongly on the second factor and only one item loaded strongly on the third. The four-factor solution did not converge. Therefore, a one-factor solution was deemed to be the most appropriate fit for the data. Inspection of the modification indices suggested that a correlated error should be added between “When I remember things we did together it feels like I am no longer the same person” and “When I remember something I did in the past; it feels like I am no longer the same person” (MI = 47.89). These correlated errors are likely due to similar wording and as such it seems likely that some of the shared variance between these items can be attributed to item wording instead of the factor on which these items load , χems load .N = 328). The fit statistics for the one-factor indicated a good fit CFI = 0.98 and TLI = 0.98, and a close to good fit for RMSEA = 0.062 (0.06–0.07), χ2 = 678.35 on df = 323, χ2:df = 2.10) supporting the chosen solution in the EFA. The CFA assessed the fit of the chosen one-factor solution with one correlated error using the CFA sample . Correlations between the total score of the OG-M and symptom measures of PGD , PTSD , depression , and behavioural proximity seeking  were all strong and significant, confirming criterion validity. The total loss-related memories scale demonstrated excellent composite reliability (ω = 0.97) and sufficient convergent validity (AVE > 0.5). Test-retest reliability over 7 days for the OG-M was excellent (N = 676), 11 items were removed from the OG-M to create the Oxford Grief–Memory Characteristics–Short Scale (OG-M-S)  for PGD ICD-11 and 0.80  for PGD DSM-5-TR which was comparable with the full OG-M ICD-11; AUC = 0.84 ; DSM; AUC = 0.80 . Converting the AUC to Cohen’s d using the formulas described in Ruscio  in the prediction of PGD  . Two of n Ruscio  gives a n of PGD .2 = 225.87 on df = 104, χ2:df = 2.17, CFI = 0.99, TLI = 0.99, RMSEA = 0.059 (0.05–0.07). The two-factor solution had a marginally better fit χ2 = 179.45 on df = 89, χ2:df = 2.01, CFI = 0.99, TLI = 0.99, RMSEA = 0.055 (0.03–0.06), but was not optimal because only two items loaded strongly on the first factor. Confirmatory factor analysis of the one factor solution resulted in a good fit CFI = 0.99, TLI = 0.99, χ2 = 200.32 on df = 104, χ2:df = 1.92, RMSEA = 0.055 (0.04--0.07)1.Exploratory factor analysis of the shortened scale revealed one eigenvalue above 1 (9.09). The one factor solution was a good fit to the data χr = 0.87. Criterion validity correlations with the symptom scales were all strong and significant . The scale also met the requirements of convergent validity AVE = 0.67.The shortened scale had excellent composite reliability ω = 0.97 and good test-re-test reliability 2 = 1.47, df = 2, p > 0.05, RMSEA = 0.00 (0.00–0.11), CFI = 1.00, TLI = 1.00; DSM, χ2 = 5.12, df = 2, p > 0.05, RMSEA = 0.075 (0.00–0.16), CFI = 1.00 TLI = 0.98). Parameter estimates are presented for PGD ICD-11 in 2.Both the PGD ICD-11 and the PGD DSM-5-T5 cross-lagged models were an excellent fit to the data . It was designed to measure the content, triggers, qualities, and consequences of loss-related memories and was developed from interviews with bereaved individuals with and without a diagnosis of PGD . ExploraAn important finding was that while early memory characteristics and PGD symptoms 0–6 months) showed significant cross-lagged relationships, each measure at baseline predicting the other 6 months later (short-term follow-up), only memory characteristics at 6–12 months predicted the long-term maintenance of PGD symptoms at 12–18 months, whereas symptom severity at 6–12 months no longer predicted memory characteristics at long-term follow-up. This pattern of results would be expected by psychological models of PGD  months s–53, 67.Factor analyses suggested that the memory characteristics measured with the OG-M are explained by one underlying dimension. This could be partly because the sample included a wide range of bereaved people with a wide range in severity of their grief reactions and endorsement of the memory items. It is possible that in a clinical sample of patients with PGD, factor analyses would reveal a more complex factor structure. However, it is also possible that a unidimensional factor structure would be retained as the items of the OG-M can be interpreted as representing different aspects of memory “re-experiencing” driven by poor loss memory integration.Previous work has suggested that traumatic memories, including traumatic loss memories, that are poorly integrated with other information in autobiographical memory are easily triggered due to poor inhibition of cue-driven retrieval. As a result, varied sensory information that has similarities with the traumatic situation has the ability to trigger intrusive trauma memories that appear to happen in the present , 11. FurAnother potential indicator of the failure to process loss memories as permanent and final measured by the OG-M is the intrusive desire to both search for and comfort the deceased . Such inThe OG-M and OG-M-S provide a comprehensive research and clinical tool for measuring aspects of loss memories relevant to poor loss memory integration. However, there are some limitations that should be considered. The study used online self-report measures to measure psychopathology making the probable diagnoses reported here less reliable than formal clinical interviews. Also previous research has suggested that self- report measures may overestimate the prevalence of mental health problems . TherefoDespite these limitations, this study presents the comprehensive OG-M and shortened OG-M-S questionnaires for the assessment of loss-related memory characteristics after bereavement and demonstrates their capacity to serve as metrics for poor loss memory integration relevant to the development and maintenance of PGD and importantly, its treatment.ethics@medsci.ox.ac.uk Readers can request analyses scripts and corresponding outputs from the Centre for Anxiety Disorders and Trauma at the University of Oxford, The Old Rectory, Paradise Square, Oxford, OX1 1TW. Email: oxcadat.enquiries@psy.ox.ac.uk.The datasets presented in this article are not readily available because no data arising from this study can be shared publicly in line with the terms of the ethics agreement provided by the University of Oxford Medical Sciences Inter-Divisional Research Ethics Committee (MS-IDREC-C1-2015-230 and MS-IDREC-C1-2015-231), which stipulated that no data could be shared with anyone outside of the research team. The relevant ethics board can be contacted on The studies involving human participants were reviewed and approved by the University of Oxford Medical Sciences Inter-Divisional Research Ethics Committee (MS-IDREC-C1-2015-230 and MS-IDREC-C1-2015-231). The patients/participants provided their written informed consent to participate in this study.KS and AE developed the study concept and design. KS performed the data collection and statistical analysis under the supervision of AE. KS drafted the manuscript. AE and JW provided the critical revisions. All authors contributed to the questionnaire development.The views expressed are those of the author(s) and not necessarily those of the NHS, the NIHR, or the Department of Health. These funding sources did not have a role in the design of the study, collection, analysis, and interpretation of the data and in writing the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


HDL subclass distributions were analyzed using the Quantimetrix Lipoprint® system. The HDL from the elderly subjects exhibited a lower level of CEC, at 11.12% (p < 0.0001), than the HDL from the young subjects. The CEC of the elderly subjects returned to normal levels following 12 weeks of EVOO intake. An analysis of the distribution of HDL subclasses showed that HDL from the elderly subjects were composed of lower levels of large HDL (L-HDL) (p < 0.03) and higher levels of small HDL (S-HDL) (p < 0.002) compared to HDL from the young subjects. A multiple linear regression analysis revealed a positive correlation between CEC and L-HDL levels (r = 0.35 and p < 0.001) as well as an inverse correlation between CEC and S-HDL levels (r = −0.27 and p < 0.01). This correlation remained significant even when several variables, including age, sex, and BMI as well as low-density lipoprotein cholesterol (LDL-C), high-density lipoprotein cholesterol (HDL-C), and glucose levels  were accounted for. Consuming EVOO for 12 weeks modulated the age-related difference in the distribution of HDL subclasses by reducing the level of S-HDL and increasing the level of intermediate-HDL/large-HDL (I-HDL/L-HDL) in the elderly subjects. The age-related alteration of the CEC of HDL was due, in part, to an alteration in the distribution of HDL subclasses. A diet enriched in EVOO improved the functionality of HDL through an increase in I-HDL/L-HDL and a decrease in S-HDL.High-density lipoproteins (HDL) maintain cholesterol homeostasis through the role they play in regulating reverse cholesterol transport (RCT), a process by which excess cholesterol is transported back to the liver for elimination. However, RCT can be altered in the presence of cardiovascular risk factors, such as aging, which contributes to the increase in the incidence of cardiovascular diseases (CVD). The present study was aimed at investigating the effect of extra virgin olive oil (EVOO) intake on the cholesterol efflux capacity (CEC) of HDL, and to elucidate on the mechanisms by which EVOO intake improves the anti-atherogenic activity of HDL. A total of 84 healthy women and men were enrolled and were distributed, according to age, into two groups: 27 young (31.81 ± 6.79 years) and 57 elderly (70.72 ± 5.6 years) subjects. The subjects in both groups were given 25 mL/d of extra virgin olive oil (EVOO) for 12 weeks. CEC was measured using J774 macrophages radiolabeled with tritiated cholesterol (( Cardiovascular disease (CVD) is one of the major causes of mortality and morbidity in developed, middle, and low income countries . The incAn increase in high-density lipoproteins (HDL) concentrations was proposed as an additional strategy to reduce the incidence of ASCVD. Epidemiological studies showed that there is an inverse correlation between plasma HDL levels and ASCVD risk . However3 (7.8–8.8 nm), and (2) large, light, and relatively cholesterol-rich particles, classified as HDL2 (8.8–12 nm) [HDL are macromolecular complexes composed of different subclasses. Several techniques were developed to distinguish or separate the different HDL subclasses according to electrophoretic mobility, size, apoprotein content, or density. The separation of HDL by density gradient ultracentrifugation makes it possible to discriminate two main subclasses: (1) small, dense, and relatively cholesterol-poor particles, classified as HDL8–12 nm) . However8–12 nm) . Nuclear8–12 nm) ,12,13. HSeveral cardioprotective properties are ascribed to HDL, including cellular cholesterol homeostasis and anti-inflammatory and antioxidant activities. The former is attributed to reverse cholesterol transport (RCT), a pivotal HDL pathway whereby excess cholesterol from peripheral cells, such as arterial wall cells, is transported back to the liver to be excreted in the bile or metabolized into bile salts before excretion . RCT conSeveral studies have shown that there is an inverse correlation between the CEC of HDL and early risk markers for subclinical atherosclerosis and the incidence of cardiovascular events ,16,17. TExtra virgin olive oil (EVOO), one of the major components of Mediterranean diet, is associated with protection against cardiovascular risk factors  and this® HDL kits were obtained from Quantimetrix . Extra virgin olive oil (EVOO) was obtained from Atlas Olive Oils sarl .Sodium dodecyl sulfate (SDS), ethylenediaminetetraacetic acid (EDTA), and polyethylene glycol (PEG) were purchased from Sigma-Aldrich . Bovine serum albumin (BSA), fetal bovine serum (FBS), Dulbecco’s modified eagle medium (DMEM), Roswell Park Memorial Institute (RPMI) 1640 medium, and penicillin/streptomycin were purchased from Wisent . J774 macrophage-like cells were obtained from the American Type Culture Collection . Bio-Rad Protein Assay kits were obtained from Bio-Rad Laboratories . LipoprintEighty-four healthy men and women, aged between 23 and 85 years, were enrolled and distributed into two groups: 28 young (23–45 years) and 56 elderly (65–85 years) healthy subjects per group. All participants had a normal blood pressure and serum lipid profile. They were all non-smokers and not taking medication, including lipid-lowering drugs or antioxidant supplements. Pre- and postmenopausal women were included and none of the women in the elderly group were taking estrogen replacement therapy for menopause. The subjects were invited to consume 25 mL/d of raw EVOO for 12 weeks. Blood tests were completed at enrollment (T0) and after 12 weeks of EVOO intake (T12). The chemical composition of EVOO used in the present study is described in The present study was conducted according to the guidelines set out in the Declaration of Helsinki. The protocol was approved by the Ethics Committee of the Sherbrooke University Institute of Geriatrics (# 2009/19). Written informed consent was obtained from all subjects.Fasting blood samples from the subjects were collected in EDTA and heparin blood collection tubes to determine low-density lipoprotein cholesterol (LDL-C), high-density lipoprotein cholesterol (HDL-C), total cholesterol , triglyceride (TG), and glucose levels at T0 and T12. The blood was centrifuged to obtain plasma, with one part of plasma used to isolate HDL for CEC measurement, while the other part was stored at −80 °C until used to analyze the distribution of HDL subclasses.g for 20 min at 4 °C and the supernatant (HDL) was then collected. Total protein concentration of HDL was determined using the Bio-Rad protein assay kit with spectroscopy (595 nm) measurement. HDL concentrations were expressed in total protein concentration.Plasma samples containing EDTA from the elderly subjects were immediately used to isolate HDL (at T0 and at T12) using the polyethylene glycol (PEG) precipitation method. The plasma and PEG solution  were incubated for 20 min at room temperature. The mixture was centrifuged at 2000× HDL subfractions were measured using a Quantimetrix Lipoprint® system HDL subfractions kit according to the manufacturer’s instructions. Briefly, 25 µL of plasma sample from each subject were placed on the upper part of individual high-resolution 3% polyacrylamide Lipoprint® HDL gel tubes. The gels were photopolymerized at room temperature for 30 min. The electrophoresis was performed using a 3 mA current per gel tube at 500 V for 55 min, after which the tubes were left to rest for 30 min. The tubes were then scanned, and the digitized images were analyzed with Lipoware software. For the determination of the amount of cholesterol (mg/dL) in each band, the HDL-C concentration of the sample was multiplied by the relative area of each HDL subfraction. This method allowed us to identify 10 HDL subfractions, which could be regrouped into three subclasses: large HDL (L-HDL), medium HDL (I-HDL), and small HDL (S-HDL) particles corresponding, respectively, to subfractions 1 to 3, subfractions 4 to 7, and subfractions 8 to 10. L-HDL and I-HDL/S-HDL were also defined as fractions with densities comprised between 1.063–1.125 g/mL and 1.125–1.21 g/mL, respectively ,23.3H) cholesterol for 24 h. The labelled cells were washed and equilibrated in a serum-free medium (1% BSA) for 12 h. They were then washed and incubated for 24 h with or without HDL (50 µg/mL) from individual subjects. At the end of the incubation period, the growth medium was recovered and centrifuged to remove cell debris. The cells were washed and thereafter lysed in PBS containing 1 M NaOH. The (3H) cholesterol levels were measured in the medium and the cellular lysates using a liquid scintillation counter . The level of cholesterol efflux (expressed in percentage) was calculated using the following formula: The cholesterol efflux measurements were based on a previously described method . Brieflyp < 0.05 was considered statistically significant.Data are expressed as mean ± standard deviations (SD). A Student’s t-test and nonparametric tests were used when appropriate to determine statistically significant differences between the young and elderly group results. An ANOVA , followed by a Tukey multiple comparisons test, was used to compare more than two groups. Pearson’s correlation coefficient was used to assess the correlation between CEC, age, S-HDL, and L-HDL after adjusting for possible confounding variables , and blood pressure, etc.). The statistical analyses were performed using GraphPad Prism v9.0 . For all analyses, p < 0.001). The elderly subjects presented slightly, but significantly higher, BMIs (p < 0.03); higher total cholesterol (p < 0.001), LDL-C (p < 0.001), and triglyceride (p < 0.03) levels; higher glycemic index; and higher systolic and diastolic blood pressures (p < 0.03) at baseline compared to the young subjects. These values were, however, within physiological ranges. EVOO consumption for 12 weeks resulted in a trend toward a lower mean for several of the parameters, including total cholesterol, LDL-C, and triglycerides as well as systolic and diastolic pressures. However, these decreases were not statistically significant  for 12 weeks. The main biochemical and anthropometric data of all the subjects are presented in nificant .p < 0.03) compared to the baseline values  compared to HDL from the young subjects (p < 0.02) after 12 weeks of EVOO consumption, whereas no significant change in the CEC of HDL from the young subjects was observed between baseline and after 12 weeks of EVOO intake (p < 0.03) when compared to that of the HDL from young subjects subjected to the same diet  and after 12 weeks (T12) of EVOO consumption for the young and elderly subjects. At baseline, HDL from the older subjects exhibited a lower capacity to mediate cholesterol efflux  (r = −0.012 and p < 0.24)  A. Howeve < 0.24) B.To elucidate the mechanisms that may explain the improvement in the CEC of HDL in response to EVOO intake, despite the absence of any change in plasma HDL concentrations, we investigated the effect of EVOO intake on the distribution of HDL subclasses. Plasma from each subject was subjected to electrophoresis in non-denaturing polyacrylamide gradient gels using the Lipoprint system. The distribution of the HDL subclasses showed that L-HDL, I-HDL, and S-HDL made up 23.9% ± 14.95%, 52.52% ± 13.10%, and 22.7% ± 18.99%, respectively, of the total HDL . No signp < 0.002) and lower levels of L-HDL (p < 0.03) compared to the young subjects (p < 0.0004).We then investigated the effect of EVOO intake as a function of the age of the participants (young vs. elderly) on the distribution of the HDL subclasses. subjects A. No difsubjects B. EVOO isubjects C. However = 0.35 and p < 0.001) and negatively and significantly correlated with S-HDL levels (r = −0.27 and p = 0.01) (r = 0.31 and p < 0.001), but was no longer significant for S-HDL levels . However, the CEC of HDL was positively and significantly correlated with L-HDL levels A (r = 0. = 0.01) B. The coL levels C,D. No rp < 0.002 and β = −0.24, p = 0.01, respectively)  .The development of atherosclerosis and the incidence of its risk factors and clinical manifestations increase dramatically with age and are responsible for most cardiovascular morbidity and mortality in the elderly. According to the Framingham risk score, the risk of developing CVD within ten years increases by one point every five years . This ma2 and an increase in HDL3 [We evaluated the functionality of HDL by measuring the capacity of HDL to mediate cholesterol efflux, which is a metric of HDL functionality and a rate-limiting step of the RCT process. Previous studies showed that CVD is associated with an alteration in the CEC of HDL  and a ch in HDL3 . Our res in HDL3 ,29. In tOur results showed that HDL from the healthy elderly subjects in our cohort had higher S-HDL and lower L-HDL levels than the young subjects. This change in the distribution of HDL subclasses could be explained by the biochemical changes that occur during aging. Numerous studies showed that HDL from elderly people were characterized by significant changes in their protein and lipid composition compared to HDL from younger people . Indeed,3 (S-HDL) levels were associated with decreased cardiovascular risk, indicating that HDL3 is cardioprotective [3 remained equivocal [2 (L-HDL) subfractions were associated with increased protection against atherosclerosis [2 and HDL3 subclasses and the stage of arterial atherosclerosis [Comparative studies of the anti-atherogenic effect of S-HDL and L-HDL have given conflicting results. An analysis of data from the Framingham study showed that plasma HDLotective . Conversquivocal  and may clerosis ,36. Lastclerosis . NeverthAlthough S-HDL initiate the unidirectional transfer of cellular cholesterol via the apoA1-ABCA1 pathway, L-HDL interact with ATP binding cassette transporter A1, ATP binding cassette sub-family G member 1 and scavenger receptor class 1  and mediate the bidirectional exchange of cellular cholesterol. Consequently, a high level of S-HDL in the elderly may be the result of a weak interaction between S-HDL and peripheral cells (apolipoprotein A1 (apoA1)-ABCA1 pathway), which reduces the maturation of S-HDL to L-HDL in vivo. This is supported by the work of Sene et al., who showed that there was an age-related reduction in ABCA1 expression, which causes an impairment of senescent macrophage cholesterol efflux . Higher Interestingly, our results showed that the CEC of HDL from the elderly subjects improved after 12 weeks of EVOO intake. Cholesterol efflux is a process involving the transfer of cellular cholesterol to HDL particles, which then transport the cholesterol to the liver for elimination. Cholesterol efflux is thus the fundamental process in the protection from the development of atherosclerotic lesions. Several studies showed that there is a clear association between lower cholesterol efflux and the risk of developing ACVD. Cholesterol efflux depends both on the ability of macrophages to liberate cholesterol and on the capacity of HDL to accept cholesterol. Although the liberation of cholesterol depends on the expression level of membrane cholesterol receptor (SR-BI) and transporters (ABCA1 and ABCG1), the CEC of HDL depends on the composition, physicochemical properties, and oxidative status of HDL . The levAlthough the present study was limited to investigating the link between the distribution of HDL subclasses and their CEC, changes in the biochemical composition and structure of HDL during the maturation process allow them to acquire other cardioprotective functions. HDL subclasses also differ in their antioxidant and anti-inflammatory activities. In vitro studies showed that L-HDL displayed significantly greater antioxidant activity than S-HDL ,43 and tr = 0.244 for CEC and HDL-L, and r = 0.273 for CEC and S-HDL) [The positive correlation between the level of L-HDL and CEC and the negative correlation between the level of S-HDL and CEC indicates that L-HDL have an anti-atherogenic effect and that S-HDL have a pro-atherogenic effect. Associations between HDL size and CEC were studied, but in different contexts. In a cross-sectional study, Mutharasan et al. showed that S-HDL and L-HDL levels were negatively and positively correlated with CEC, respectively . The samd S-HDL) . This sud S-HDL) , that a 3, and HDL2), unlike other studies that, due to the separation of HDL by ultracentrifugation, excluded the pre-β HDL fraction [Our results showed that EVOO intake significantly improved the CEC of HDL in the elderly, bringing it to levels similar to that measured for HDL from young people. In addition, EVOO intake induced a change in the distribution of HDL, significantly reducing S-HDL and increasing I-HDL levels in the elderly. The level of L-LDL also increased after EVOO intake; however, this increase was not statistically significant. Our results are in agreement with those of Hernáez et al., who used randomized clinical trials to show that consuming 25 mL/day of EVOO improved the CEC of HDL ,48. Howefraction .The improvement in CEC by HDL following EVOO intake could be explained by the physicochemical changes that occur within HDL particles following EVOO intake. Indeed, several studies showed that EVOO has a beneficial effect on various properties of HDL associated with functionality, including chemical composition and the activities of HDL-associated enzymes ,48. ThesEVOO intake is also associated with a more stable conformation of apoA-1, the main component of HDL particles, and may help explain the improvement in CEC . We showEVOO intake improves the distribution of HDL subclasses, particularly in the elderly, by increasing the level of I-HDL and L-HDL and decreasing the level of S-HDL. The beneficial effect of EVOO on HDL maturation can be explained by different mechanisms, more particularly by its effect on the activity of enzymes associated with HDL, including lecithin-cholesterol acyltransferase (LCAT) and cholesteryl ester transfer protein (CETP). LCAT plays an important role in the maturation of HDL by promoting the esterification of free cholesterol, gradually converting nascent HDL (S-HDL) into mature HDL (L-HDL) . HernáezCholesteryl ester transfer protein (CETP) facilitates the transfer of cholesterol from HDL to VLDL/LDL , which pIn summary, our results showed that there was a significant decrease in the CEC of the HDL of elderly healthy subjects. Given the importance of cholesterol efflux as the first step in the RCT process, a decrease in the CEC of HDL can have a significant impact on cellular cholesterol homeostasis and the development of the atherosclerotic process in the elderly. We also showed that the distribution profile of HDL subclasses in the elderly, even if healthy, was altered to a more pro-atherogenic HDL distribution profile (decrease in L-HDL and increase in S-HDL). Lastly, our results showed that a short 12-week intervention with an EVOO-rich diet stimulated the CEC of HDL to a level comparable to that of young, healthy subjects and improved the distribution of HDL subclasses.Our study presents two limitations: (1) although, the participants were invited to maintain the same lifestyle  throughout their participation to the study, it would have been very helpful to evaluate the nutritional differences between participants and, particularly, their intake of olive oil, regardless of the supplemented one, and to measure the levels of physical activity for each participant; and (2) the number of young people was very low when compared to elderly people and this may explain why some results were at the limit for significance. An increase in the number of participants in the young group would have allowed us to also make comparisons in each group according to sex.


Purpose: This study evaluated Japanese patients' adherence to first-line therapy and physicians' compliance with the guidelines (GLs) for glaucoma in daily practice.Methods: This retrospective, noninterventional cohort study used a Japanese health care/pharmacy claims database from 2011 to 2016. We compared adherence based on the primary outcomes assessed as proportion of days covered (PDC) and persistence between patients who received first-line monotherapy followed by second-line fixed-combination therapy (GL-compliant cohort) and those who received first-line fixed-combination therapy (GL-noncompliant cohort). Furthermore, we explored treatment patterns, glaucoma consultation, and key factors associated with GL noncompliance.Results: Among 11,666 patients who received first-line therapy, 1,183 (10.1%) and 542 (4.6%) were in the GL-compliant and GL-noncompliant cohorts, respectively. Prostaglandin (70.7%) and subsequent prostaglandin/β blocker (BB) (20.9%) within 12 months were most used by the GL-compliant cohort. Prostaglandin/BB (48.9%) and carbonic anhydrase inhibitor/BB (51.1%) were prescribed for the GL-noncompliant cohort. The mean PDC [standard deviation (SD)], persistence rate , and consultation (SD) over 12 months were 60.9% (34.0), 16.0% (14.0–18.1), and 5.23 (3.21) times, respectively, in the GL-compliant cohort and 59.7% (35.8), 22.0% (18.6–25.5), and 4.76 (3.19) times, respectively, in the GL-noncompliant cohort. No significant differences were observed between the 2 cohorts. No clinically relevant factor associated with GL noncompliance was found.Conclusions: Around 5% of patients were prescribed a fixed-combination eye drop as first-line therapy not in accordance with GLs. The similarity of adherence and persistence between the 2 cohorts indicates that first-line fixed-combination therapy could be considered for glaucoma treatment. The proportion of patients with blindness caused by glaucoma increased to 28.6% in the fiscal year 2015 from previous surveys in the fiscal years 2007–2009.4 However, some patients may be better to start receiving multidrug or laser treatment since it is also known that the disease may continue to progress during the late stages of glaucoma, even when treatment is provided. Accordingly, early detection and treatment of glaucoma are crucial. A recent study reported that selective laser trabeculoplasty was more cost-effective than eye drop treatment as a first-line therapy,5 although the medical costs vary from country to country.The Japan Glaucoma Society Guidelines for Glaucoma (4th edition) state that reduction of intraocular pressure (IOP) is the only established treatment for primary open-angle glaucoma (broad definition of glaucoma), and the guidelines (GLs) recommend starting with a single drug (monotherapy) as first-line therapy, 2 drugs as second-line therapy, increasing by one drug each time, followed by laser treatment or surgery.6 Another randomized clinical trial demonstrated that fixed-combination therapy (travoprost/timolol) improved long-term adherence of patients with glaucoma compared with unfixed therapy (travoprost and timolol).7 Our database study also demonstrated that fixed-combination therapy was associated with greater adherence than unfixed therapy.8Adherence to eye drops is clinically critical and is influenced by the number of eye drops used in daily practice. A previous 6-month observational trial reported that a fixed-combination (latanoprost/timolol) therapy led to a superior adherence rate and ocular surface health compared with unfixed-combination (latanoprost and timolol) therapy.9 and was developed with the aim of reducing the number of eye drops (reducing patient burden), avoiding the washout effect due to an inadequate interval of instillation time (reducing the drug's efficacy), and reducing preservative exposure (reducing the risk of ocular surface disease). As such, fixed-combination eye drops have advantages in glaucoma treatment and can be used as a first-line therapy at the physician's discretion in daily practice. However, since the release of the GLs, there have been no large-scale studies since in Japan on the state of GL compliance in clinical practice or adherence to first-line treatment.Specifically, latanoprost/timolol fixed-combination therapy for glaucoma has a high adherence rate. It lowers IOP more potently than unfixed-combination latanoprost or timolol monotherapy,Our primary objective was to compare patient adherence and persistence with first-line therapy between GL-compliant and GL-noncompliant cohorts using a Japanese administrative health care/pharmacy claims database. Secondary objectives were to explore physicians' compliance with the GLs for glaucoma, the pattern of annual glaucoma treatment, incidence of annual glaucoma consultation, incidence of annual glaucoma surgery, and key factors associated with GL noncompliance.This retrospective, observational cohort study used a commercially available Japanese health care/pharmacy claims database to investigate the prescription status in daily practice. The database did not include data for glaucoma severity  that may influence the treatment course of glaucoma (monotherapy or fixed-combination therapy as a first-line therapy). To address this limitation, we designed the distribution of glaucoma severity to be as comparable as possible between the cohorts. Therefore, data for patients with glaucoma who were newly prescribed a topical IOP lowering medication as a first-line therapy of either monotherapy followed by second-line fixed-combination therapy (GL-compliant cohort) or first-line fixed-combination therapy (GL-noncompliant cohort) were extracted and compared between the cohorts.10 set out by the Japanese government. This research project adhered to the tenets of the Declaration of Helsinki.Since this study used only anonymized data from patients and did not use information linkable to individual identification, ethical approval and informed consent from the patients were not required according to the “Ethical Guidelines for Medical and Health Research Involving Human Subjects”12 The database is regularly updated and has accumulated data since 2008 and, as of April 2017, included data from 6.3 million individuals, accounting for 1.7% of the Japanese population.13 Furthermore, it is generally consistent with 2 other national databases, and is useful as it has a low selection bias and a large sample size with a wide age distribution; however, it only targets large nationwide corporations and does not include individuals in primary industries or the self-employed.11The database was provided by MinaCare Co., Ltd., and secondary use of the data was permitted by each health insurance group. The MinaCare database contains anonymized data on both health checkup and medical/pharmacy claims of workers and their family members among a wide range of age groups <75 years.Patients with International Classification of Diseases, 10th Edition (ICD-10) diagnosis codes H401  and H409 (unspecified glaucoma) were extracted from the database between April 1, 2011 and March 31, 2016. Patient identification (anonymized), age, sex, body weight, height, body mass index (BMI), smoking status, ICD-10 diagnosis codes, name/date of any prescribed drugs, hospital location, name of glaucoma medication class, name/date of glaucoma diagnosis tests, name/date of glaucoma-related surgery, and health checkup data, such as anthropometric measurements, blood pressure, blood glucose level, blood lipid level, liver function test values, hematologic values, and urine test results from the health checkup database, were also extracted. The index date was defined as the date of first prescription of topical IOP lowering medication for each patient, and the study was divided into 2 periods, before and after the index date, to determine newly prescribed medication for each patient. Among the extracted patients with glaucoma, eligible patients included those who did not receive glaucoma treatment in the preindex period for ∼12 months, and thereafter had at least one record of prescription of topical glaucoma medication on or after the index date for ≥1 year. Topical glaucoma treatments included prostaglandin (PG) analog, β blocker (BB), carbonic anhydrase inhibitor (CAI), α2 receptor agonist (AA), α1 receptor blocker (AB), αβ blocker (ABB), cholinergic agonist, muscarinic agonist (MA), rho-associated coiled kinase inhibitor (ROCKI), fixed combinations and their generics. Patients with a history of glaucoma surgery, including laser treatment, on or before the index date were excluded.The GL-compliant cohort was defined as patients who were newly prescribed a monotherapy as first-line therapy followed by a fixed-combination therapy as second-line therapy during the postindex period. Patients who switched to an unfixed-combination therapy were excluded. Patients who switched to other drug class monotherapies were considered adherent to first-line monotherapy. Patients who switched to a combination therapy were considered to have discontinued their treatment to start second-line therapy, indicating that they required more potent or appropriate medication. In contrast, the GL-noncompliant cohort included patients with a fixed-combination therapy as first-line therapy. Patients who switched to an unfixed-combination therapy or monotherapy were considered to have discontinued their treatment.a priori definition of the adherence cutoff point was 80%. Secondary outcomes were physicians' compliance with the GLs for glaucoma, the pattern of annual glaucoma treatment, incidence of annual glaucoma consultation, and incidence of annual glaucoma surgery. The key factors associated with noncompliance to local GLs were also explored using patients' demographic and clinical characteristics at the index date.The primary outcomes of this study were adherence and persistence rates with the first-line topical glaucoma medication over the 12-month postindex period. Adherence was measured by the proportion of days covered (PDC) with medication. The 14 2 areas (East Japan vs. West Japan),15 and city size (big cities including Tokyo and 20 ordinance-designated cities compared with the others).16Patient background characteristics included sex, age, body weight, height, BMI, smoking status/history, health checkup results, comorbidity, and residency region. Residency regions were classified into 8 regions from Hokkaido to Kyushu, including Okinawa,PDC was calculated for first-line topical glaucoma medication (monotherapy or fixed-combination therapy), regardless of switching to other drug classes. To address possible variations in dosage and the prescribed period reported on a claim for glaucoma medication, adjustments were made based on the unit volume of the formulation. For example, if the dosage was “7.5” and the period was “1” in a claim, this was converted to “2.5 mL (approved bottle) × 3 bottles” prescribed for 3 months, since one bottle (2.5 mL for once-daily formulations or 5 mL twice-or-more-daily formulations) is sufficient for application to one eye for >30 days but corresponds to 30 days dispensed. In Japan, the recommended use by period of an opened bottle of eye drop is generally 4 weeks. Accordingly, patients were permitted a 30-day grace period to obtain the next prescription. In addition, if one prescription consisted of ≥7 bottles, the patient was considered as receiving treatment for both eyes. The PDC value was calculated as follows:Persistence refers to the act of continuing index therapy . The definition of “persistence” is the duration in days from the index date (the first prescription date of the index therapy) to the last prescription date plus the number of prescription days or to the discontinuation date allowing a 30-day grace period. Treatment was considered as “discontinued” if there was no prescription record for >30 days or if the patient switched to second-line prescriptions. The cumulative discontinuation rate was analyzed based on the time to discontinuation and was estimated using the Kaplan–Meier method. Patients who continued the first-line treatment for 12 months were treated as censored at 12 months. The persistence rate at the evaluated time point was calculated as follows:The annual consultation rate was calculated as the number of eye examinations per patient for 12 months after the index date. The examinations were categorized into 3 groups: (1) general eye examination including ophthalmoscopy  and tonometry (IOP test); (2) perimetry ; and (3) optical coherence tomography.t test. In addition, the number and proportion (percentages) of patients in the following categories were identified by cohort: PDC ≥0.8; PDC <0.8; 0.6 ≤ PDC <0.8; 0.4 ≤ PDC <0.6; 0.2 ≤ PDC <0.4; and PDC <0.2. The proportion of patients with or without PDC ≥0.8 between the 2 cohorts was compared using Pearson's χ2 test.Results were presented as mean and standard deviation (SD) or number and proportion of patients. The mean PDC for 12 months between 2 cohorts was compared using Time (in days) to treatment switching or discontinuation over 12 months was analyzed using the Kaplan–Meier method, and the persistence rate at 6 and 12 months and the corresponding Wald 95% confidence intervals (CIs) were calculated. In addition, differences in persistence were compared using the log-rank test. Patients who did not switch to other medications or did not discontinue first-line medication were censored on the last day of the 12th month in the postindex period.m = 6, where m is the total number of statistical tests in this study). All P values >0.05/6 ( = 0.00833) were considered statistically significant for comparisons between cohorts.Bonferroni correction was used to adjust for multiplicity .Univariate logistic regression model was performed with subsequent multivariate analyses for those with A total of 57,899 patients were diagnosed with glaucoma in the database . After eThe baseline demographics and clinical characteristics of the 2 cohorts were similar . The meaThe eye drops prescribed at the index date are summarized in P = 0.0061), although this did not reach statistical significance for the mean value . The persistence rates (95% CI) at 6 and 12 months postindex were 37.1% (34.4–39.9) and 16.0% (14.0–18.1), respectively, in the GL-compliant cohort, and 36.7% (32.7–40.8) and 22.0% (18.6–25.5%), respectively, in the GL-noncompliant cohort. Thus, the persistence rate over a longer period showed a better trend for the GL-noncompliant cohort. Compared with the subset of patients switching from latanoprost to latanoprost/timolol in the GL-compliant cohort, the latanoprost/timolol subset in the GL-noncompliant cohort showed a tendency toward higher persistence, but this did not reach statistical significance  in the GL-compliant cohort and 10 patients (1.8%) in the GL-noncompliant cohort, and the number of surgery cases was 0.42 and 0.71/100 person-years of observation (PYO), respectively. There were no marked differences in characteristics between 2 cohorts .P = 0.0205), and asthma or chronic obstructive pulmonary disease (P = 0.0921)  ( 0.0921) . Multivampliance . HoweverTo date, there have been no studies on GL compliance of physicians in new patients with glaucoma in Japan or other countries. This study revealed that most Japanese patients with glaucoma were treated in accordance with the GLs, but ∼5% of patients were prescribed a fixed-combination eye drop as first-line therapy. Although adherence to the therapy was moderate, there were no significant differences between the GL-compliant cohort and the GL-noncompliant cohort regarding the adherence and persistence to first-line therapy, annual glaucoma consultation, or annual glaucoma surgery. Therefore, first-line fixed-combination eye drops may be considered to be a convenient treatment option for glaucoma patients according to the patients' medical needs, especially for those who are already at the severe glaucoma stage at the initial glaucoma consultation and need early IOP lowering treatment, those who are relatively young and for whom it is important to delay blindness, or those who need personalized medication due to their lifestyle.According to the recommendation found in GLs as first-line treatment, PGs, such as latanoprost, tafluprost, or travoprost, were commonly prescribed. The most common treatment pattern in the GL-compliant cohort over the 12-month postindex period was PG (33.1%) followed by PG switching to PG/BB (20.9%). However, few patients switched to other single drugs or add-on therapies. On the contrary, >70% of the GL-noncompliant cohort received CAI/BB (38.7%) and PG/BB (36.0%) over the 12-month postindex period.17 One possible explanation for these differences in the persistence rates is the differences in the definition of “discontinued.” In the previous study, the change of the initially prescribed glaucoma eye drops was not defined as “discontinued,”17 but it was defined so in this study. In the GL-compliant cohort, patients who switched to a fixed-combination therapy from a monotherapy as first-line therapy were considered to have discontinued their treatment. In the GL-noncompliant cohort, patients who switched to an unfixed-combination therapy or monotherapy from a fixed-combination therapy as first-line therapy were considered to have discontinued their treatment. Differences in the first dosing regimen  may be another reason for the differences among the studies' results. In the previous study, 94.1% of new patients with glaucoma were medicated with single glaucoma eye drops17; however, all the patients were medicated with a single (the GL-compliant cohort) or fixed combination (the GL-noncompliant cohort) in this study. Therefore, both studies evaluated different things; the previous study evaluated the persistence rate to total glaucoma medication use, whereas this study evaluated the persistence rate to the first-line glaucoma medication use. A systematic review summarized the results from 14 published studies focusing on the persistence with glaucoma medication use, and reported that only 31% of new patients with glaucoma remained persistent 12 months after starting therapy.18 Taken together, it is likely that patients in this study included those requiring a change in treatment plan due to relatively severe glaucoma stage at the index date or those who required early IOP lowering treatment.The 12-month adherence to the first treatment was moderate in both cohorts and was similar between the 2 cohorts; the mean PDC was 60.9% in the GL-compliant cohort and 59.7% in the GL-noncompliant cohort at 12 months. Persistence over the 12-month postindex period did not differ, and the persistence rate of the first-line treatment was 16.0% in the GL-compliant cohort and 22.0% in the GL-noncompliant cohort. The persistence rates observed in this study were lower than those reported by a previous study (60.9%) in newly diagnosed Japanese patients.Patients in both cohorts frequently received glaucoma consultations; the mean number of annual consultations with ophthalmoscopy and tonometry was 5.23 times in the GL-compliant cohort and 4.76 times in the GL-noncompliant cohort. A similar tendency was observed for other examinations.Glaucoma surgery was observed to be performed at a certain rate in the GL-noncompliant cohort (0.71/100 PYO) and the GL-compliant cohort (0.42/100 PYO) for the 12-month postindex period. These surgery rates are likely due to the fact that patients who required early treatment but also required consideration of ocular allergy to eye drops, who had difficulty visiting the hospital, and who had cataract surgery were started with a fixed-combination drug and then immediately underwent surgery when they experienced difficulty in achieving reduction in IOP to prevent damage to the optic nerve.19No key factor was identified as being associated with GL noncompliance. This suggests that patients may be treated with either first-line monotherapy or fixed-combination therapy. Furthermore, physicians can consider prescribing fixed-combination agents according to the patient's convenience depending on their characteristics, such as lifestyle and medication adherence. However, this study did not consider severity of glaucoma as these data were not available in the database. Although we compared data between newly diagnosed patients who were prescribed first-line monotherapy and those prescribed a fixed-combination therapy, it remains unclear whether there was a comparable distribution of severity of glaucoma among the 2 cohorts. Physicians may decide on a prescription according to the severity of the disease; however, a fixed-combination therapy would be appropriate since dual therapy is more effective than monotherapy.In conclusion, the patients' adherence and persistence to first-line therapy were not very high, but no significant differences were found between the GL-noncompliant cohort and the GL-compliant cohort. About 5% among all the physicians prescribed a fixed dose combination as first-line therapy, which did not comply with the GL. Furthermore, the annual incidence of glaucoma consultation or glaucoma surgery was similar. No factors associated with GL noncompliance were noted among newly diagnosed Japanese patients with glaucoma. Thus, fixed-combination eye drops could be used as first-line therapy as well as second-line therapy for glaucoma. A first-line fixed-combination option would be useful for personalized medication in view of severity of glaucoma in addition to improvement of adherence and health care by understanding patient preferences in daily practice.


Model-based small area estimation methods can help generate parameter estimates at the district level, where planned population survey sample sizes are not large enough to support direct estimates of HIV prevalence with adequate precision. We computed district-level HIV prevalence estimates and their 95% confidence intervals for districts in Uganda.Our analysis used direct survey and model-based estimation methods, including Fay-Herriot (area-level) and Battese-Harter-Fuller (unit-level) small area models. We used regression analysis to assess for consistency in estimating HIV prevalence. We use a ratio analysis of the mean square error and the coefficient of variation of the estimates to evaluate precision. The models were applied to Uganda Population-Based HIV Impact Assessment 2016/2017 data with auxiliary information from the 2016 Lot Quality Assurance Sampling survey and antenatal care data from district health information system datasets for unit-level and area-level models, respectively.β1 = 0.66, r2 = 0.862), and correlation between area-level model and direct survey estimates was . The error associated with the estimates decreased by 37.5% and 33.1% for the unit-level and area-level models, respectively, compared to the direct survey estimates.Estimates from the model-based and the direct survey methods were similar. However, direct survey estimates were unstable compared with the model-based estimates. Area-level model estimates were more stable than unit-level model estimates. The correlation between unit-level and direct survey estimates was (Although the unit-level model estimates were less precise than the area-level model estimates, they were highly correlated with the direct survey estimates and had less standard error associated with estimates than the area-level model. Unit-level models provide more accurate and reliable data to support local decision-making when unit-level auxiliary information is available. Model-based small area estimation (SAE) methods can help monitor the impact of public health interventions and appropriately allocate resources in small geographical areas where the domain-specific sample is not large enough to support direct estimates of adequate precision. Other terms used to refer to small geographical areas include “small domain” or “local area” . SAE metSources of auxiliary information may include routine administrative data from Health Information Systems (HMIS), censuses or other surveys. However, external covariate data are limited for predicting local area estimates. For example, routine data only capture information from individuals who interact with health facilities –4. A stuThe LQAS methodology is recommended as a tool for monitoring public health interventions , 14. In Model-based SAE methods are classified into two types: 1) Unit-level models such as the Battese-Harter-Fuller model , which lSAE studies in Africa have been limited to area-based models to estimate institutional births in Ghana ; to idenUganda’s HIV prevalence distribution varies across geographical regions and among population groups. National HIV prevalence is 6.2% among persons aged 15–64 years  and varies from 3.1% to 8.0% across regions . AvailabWe applied unit-level and area-level models to estimate HIV prevalence for districts in Uganda. We compared the precision of the model-based estimates to direct survey estimates and the precision of the unit-level model to the area-level model estimates. Our findings include several alternative district-level estimates that may be helpful for monitoring district-level HIV/AIDS intervention programs.The Uganda Population HIV Impact Assessment (UPHIA) 2016–2017 used a two-stage, stratified cluster-sampling design. In the first stage, 520 enumeration areas (EAs) or clusters were randomly selected using probability proportional to the number of households in the cluster; in the second stage, a sample of 25 households were randomly selected using equal probabilities from each EA. The EAs were based on the 2014 National Population and Housing Census (NPHC) . For ourIn Uganda, annual LQAS surveys are used to obtain district-level indicator estimates for monitoring health interventions . In LQASSummary district level covariate data were obtained from the NPHC 2014 , and HIVWe computed district-level HIV prevalence estimates for 70 districts that conducted both UPHIA and LQAS surveys in 2016 using direct survey, area-level SAE models, and unit-level SAE models. We further assessed the estimates for consistency in estimating HIV prevalence and compared the precision of the estimates via the confidence intervals and the coefficient of variation of the estimates.yij is the HIV status  of the j-th individual in the i-th district and pi is the proportion of HIV-positive people in district i, then taking into account the sampling weights, the direct estimate of district HIV prevalence is obtained as follows:i = 1,2, …,m, j = 1,2, …,Ni.If m is the number of districts and Ni is the number of individuals in district i.Where Using the direct survey estimate , we usedThe multivariate Fay-Herriot model  is the m2.1. Model estimation and prediction. Taking p area-level covariates, zi, where zi =  for the m areas, the linking model for the area-level parameter θi is defined asβ = T is a p × 1 vector of regression coefficients and E(υi) = 0 and υi, capture the unstructured heterogeneity among the areas (districts) not explained by the sampling error variance.θi is obtained using a sampling model in model 2.2.i = 1,2, …,m, ei ~ N. Where ei is the sampling error with known sampling variance, Var (ei) = ψi and E(ei) = 0 for all areas.The unbiased direct estimator of θi is unobservable and is estimated based on the sampled observations in the area.Model 2.2 is referred to as the sampling model because ei ~ N, i = 1,2,…,m and υi is independent of eiCombining 2.1 and 2.2, we obtain the mixed modelThe Fay-Herriot, area-level model estimate is then obtained as a weighted combination of the direct  was found to be related to HIV prevalence from ANC attendance (PANCi) yi = logit(pi) and zi = logit(pANCi). This model had the lowest value of the Akaike Information Criterion (AIC)  conditioned on pij’s with area random effects υi; β is the vector of regression model parameters. υi is assumed to be independent and identically distributed with E(υi) = 0 and imate pi , 32.loAssuming absence of area-level auxiliary information, the indirect estimators of district HIV prevalence, based on model 3.1 is the empirical best predictor obtained as follows:ni values of HIV infection for sampled individuals from the i-th district while i, and Ni corresponds to number of individuals in each district. Model fitting and parameter estimation were implemented using the SAE [The component  the SAE  package  the SAE The MSE of ulations , 34 as d3.2. Auxiliary variables for the unit-level model. Auxiliary variables from the 2016 LQAS data include age group in years , sex , level of education , marital status  and number of sexual partners including spouse in the 12 months preceding the survey . We selected these variables because they were significantly associated with HIV positivity in our previous study [We used summary statistics, regression analysis, and graphical assessment for consistency to compare estimates from direct, area-level, and unit-level models. We assessed gain in precision of the model-based estimates compared to the direct survey estimates using the coefficient of variation of the estimate and ratio of the MSE. We further computed ratios of the unit-level model and the area-level model estimates to assess for improvement in precision of the unit-level model estimates.Ethical clearance to conduct this study was obtained from the University of Witwatersrand Human Research Ethics Committee (HREC), clearance Certificate number M171053. Further clearance was obtained from the Uganda National Council for Science and Technology (UNCST) with registration number HS2366. Data for the study were obtained from surveys conducted in Uganda. The study was a secondary analysis of data, so consent to participate is not applicable.For both surveys, most respondents were women, had incomplete primary education, were married/cohabiting, and had one sexual partner in the 12 months preceding the surveys .HIV prevalence estimates for the districts included in the analysis are summarized in Unit-level model HIV prevalence estimates β1 = 0.66; r2 = 0.862; β1 = 0.44; r2 = 0.698; We regressed model-based estimates against direct survey estimates to assess bias and reliability of the model-based estimates. The bias diagnostics plot also presents a scatter plot of the estimates and shows the effect of extreme values in the estimates. Unit-level model estimates were strongly correlated with the direct survey estimates  as observed in the national level survey . These dAlthough the model-based methods assume independence of area random effects, independence is unlikely to hold between neighboring districts. Spatial estimation approaches attempt to solve this, although these methods have their own limitations . BordersData-driven district-level decision making for HIV programs requires precise and reliable estimates, but survey sample sizes from population surveys are significantly smaller for districts than for regions. Our study shows that with external auxiliary data, estimates that are more precise can be obtained, but precision depends on whether the information is available at the sampling unit level or area level. Unit-level model estimates, although less precise than area-level estimates, were more consistent with direct survey estimates. This application also promotes use of annual community LQAS data, which is readily available to district service managers. SAE models also can be developed in freely available software, such as R and STATA. Models that use both area-level and unit-level covariates to obtain SAE could help provide more precise and accurate HIV prevalence estimates in other settings.S1 File(DOCX)Click here for additional data file.S2 File(DOCX)Click here for additional data file.S1 TableAbbreviations: CI, confidence interval.(DOCX)Click here for additional data file.


The outcome was measured using the German Pain Questionnaire (GPQ), Depression, Anxiety and Stress Scale (DASS), and SF-12 questionnaire. Pain perception significantly decreased in the DLPFC-group (38.17%) compared to the M1-group (56.11%) (p ≤ 0.001) on the later sessions. Health-related quality of life also improved in the DLPFC-group (40.47) compared to the Sham-group (35.06) (p = 0.016), and mental composite summary (p = 0.001) in the DLPFC-group (49.12) compared to M1-group (39.46). Stimulation of the left DLPFC resulted in pain relief, while M1 stimulation was not effective. Nonetheless, further studies are needed to identify optimal cortical target sites and stimulation parameters.Repetitive transcranial stimulation (rTMS) has been shown to produce an analgesic effect and therefore has a potential for treating chronic refractory pain. However, previous studies used various stimulation parameters , and the best stimulation protocol is not yet identified. The present study investigated the effects of multi-session 20 Hz (2000 pulses) and 5 Hz (1800 pulses) rTMS stimulation of left motor cortex (M1-group) and left dorsolateral prefrontal cortex (DLPFC-group), respectively. The M1-group ( Pain is recently redefined by the International Association for the Study of Pain (IASP) as “an unpleasant sensory and emotional experience associated with, or resembling that associated with, actual or potential tissue damage” . Pain isLBP with unidentifiable pathoanatomical and pathophysiological causes is the most common form of chronic pain condition and is termed non-specific low back pain (NSLBP) . With noThe most challenging issue in managing chronic pain, including NSLBP, is that the underlying mechanisms are still poorly understood. In NSLBP, pain sensations do not necessarily reflect the presence of a peripheral noxious stimulus because neurons in the pain pathway can be activated by a low threshold, innocuous or non-noxious inputs . The neuIn contrast to SI, the evidence of altered structural, organizational, and functional alternation in the primary motor cortex (M1) for neuropathic and non-neuropathic pain conditions is conflicting . HoweverIn the past decades, the development of non-invasive brain stimulation methods such as TMS and several variants of transcranial electrical stimulation allows the identification of the causal role of different cortical regions and neuromodulation of these structures to treat pathological conditions such as chronic pain. Application of repetitive magnetic pulses at a specific frequency (repetitive TMS or rTMS) or in a burst of 3–5 pulses delivered at theta frequencies (theta burst stimulation or TBS) can modulate cortical excitability during and beyond the period of stimulation ,29. InduThis study was undertaken to explore the efficacy of rTMS for NSLBP and improve the available protocol in managing chronic pain. We aimed to replicate the beneficial effect of multi-session left M1 rTMS in the study of Ambriz-Tututi and colleagues (2016). In addition to M1, we also applied neuronavigated rTMS (nrTMS) on the left dorsolateral prefrontal cortex (DLPFC) because stimulation of this brain area is reported to change pain perception in healthy subjects and has analgesic effects in acute postoperative pain, fibromyalgia, and traumatic spinal cord injury patients ,38,39. Td = 0.20, power = 0.95, and a = 0.05. Inclusion criteria were age between 18 and 80 years, clinical diagnosis of chronic LBP and or neck pain, average resting pain-level greater than 3 in the Numeric Rating Scale (0–10), no changes in pain medication 4 weeks before baseline measurements, and no single or multiple surgical procedures in the head and lower back in the last two years. Patients with the following characteristics were excluded from the study: metallic and electronic implants in the head, neck and chest; intake of opioid analgesics , tetracyclic antidepressants, antiviral, and antipsychotic drugs; history of frequent headache or tinnitus and alcohol or drug abuse; confirmed or suspected pregnancy and breastfeeding. All participants provided written informed consent before the experimental procedures. The Ethics Committee of the Medical University of Graz approved the study (registration number: 30-459-ex 17/18), and all procedures conform to the Declaration of Helsinki regarding human experimentation.Thirty-four chronic pain patients participated in the study . They were either previous neurosurgical patients or regular pain clinic patients at the University Hospital Graz-Austria. All have no prior knowledge about TMS and had no planned pain-related interventions during the study. The sample size was a priori calculated using G *Power 3.1.9 and is based on a planned repeated measure ANOVA (with within-between interaction) on the numerical pain scale data. We expected an effect size of clinicaltrails.gov (registration number: NCT04934150). The experiments took place in the outpatient clinic of the Department of Neurosurgery  between February 2019 and March 2020. The patients were allocated into an “M1-group” , “DLPFC-group” , and “sham-group”  using permuted block randomization on the online software random.org. Patients were blinded to their assigned group. Each patient in the M1-group and DLPFC-group underwent 13 nrTMS experimental sessions. The first 5 sessions were conducted every day for 5 consecutive days without a break (1 session per day). One week later, the remaining 7 sessions were conducted in a span of 9 months . The sham-group followed the same schedule; however, the experiment was stopped after the seventh session (4th week) because of ethical considerations. Each session started with head/brain and TMS coil co-registration. Subsequently, stimulation intensity was determined, and target areas underwent stimulation. Pain assessments before and after stimulation using numerical pain rating scales (NPRS) were conducted on the first (baseline), 7th (4th week), and 13th (36th week) experimental sessions  connected to a MagPro X100 stimulator . Patients were seated on a reclining chair with head and neck support and were asked to relax. For precise coil placement and stimulation, neuronavigation (line-navigated) was performed using Localite TMS Navigator software  that tracts the coil movement with an infrared stereo-optical tracking camera . The tracking system monitors the location of passive marker spheres attached to the TMS coil and head in real-time. Each patient’s T1-weighted MRI scan  was used for head and coil registration, target planning, and neuronavigation during stimulation . TMS parameters were consistent with Ambriz-Tututi et al. (2016): 2000 biphasic pulses at an intensity of 95% resting motor threshold (RMT) applied ) for 10 s at 20 Hz. RMT was determined by electromyographic recording over the abductor pollicis brevis muscle and defined as the minimum stimulator output that elicits a 50 uV motor-evoked potential (MEPs) in 5 out of 10 single-pulse TMS stimulation of M1 at rest. Anatomically defined targets over the left M1 were marked by a 5 × 2 grid overlay with 10 mm between-target spacing A, while The primary outcome variable was derived from the German Pain Questionnaire (GPQ). At baseline and for each experimental session, patients were asked to verbally rate their perceived greatest pain intensity before stimulation and bearable pain intensity after stimulation using the GPQ-NPRS ranging from 0 (“no pain”) to 10 (“worst pain imaginable”) . As secoStatistical analyses were performed using SPSS version 26 software , and figures were generated using Graphpad Prism . All data were analysed using linear mixed-effects modelling (LMM) with random intercept. We decided to use LMMs for various statistical reasons. First, LMMs are well suited for unbalanced datasets such as ours  ,46. Secot-test, we first compared the three groups  average AUC for the first five days of stimulation. Next, we compared the average AUC from the 4th to the 36th week of stimulation of the M1 and DLPFC-group.For the analysis, the GPQ-NPRS, DASS, and SF-12 scores served as the dependent variables. We first analysed GPQ-NPRS data from time points similar to when the DASS and SF-12 scores were obtained . Before the analysis, the ordinal GPQ-NPRS scores were converted into percentages of the 11-point scale . The model contained the between-subject factor “group”  and the within-subject factor “session”  and “time” (“before stimulation” = percentage of greatest perceived pain and “after stimulation” = percentage of bearable pain) as fixed factors. This analysis does not give us a complete picture of how multi-session stimulation affected pain intensity; therefore, we plotted baseline and post-stimulation GPQ-NPRS scores from each experimental session and calculated the area under the curve (AUC) using a trapezoidal method in Microsoft Excel (2019). The AUC indicates the summation of pain relief over time. Smaller AUC values indicate a greater decrease in pain intensity. Each patient’s AUC was determined, and the group average was obtained. Using an independent paired d as a measure of effect size . Significant findings from the models were explored with post hoc comparisons . Lastly, we tested collinearity in the final models by determining the tolerance and variance inflation factors. A p-value of <0.05 was considered significant for all statistical analyses. All values are expressed as the mean ± standard error of the mean (SEM).The DASS and SF-12 scores were treated as continuous variables and therefore not converted. For the DASS and SF-12 scores, the model does not contain the within-subject factor “time” since patients only answered these questionnaires after stimulation. Each patient was specified as a random factor (random intercept model) in all models. Normality (data distribution) and homogeneity of variance test were conducted using Shapiro–Wilk and Levene’s test, respectively. In cases of normality violation, all modelled data underwent logarithmic transformation (log10). We also performed a Pearson Chi-Square test to assess the model’s goodness of fit to the data. We calculated Cohen’s p ≤ 0.001), DASS scores  and SF-12 scores  are not normally distributed, hence all were log-transformed. In all models, Levene’s test showed that variances were equal for each group . Tolerance range and variance inflation factors were equal to 1.000 in the final models indicating that multicollinearity did not affect the findings.In total, 34 patients were enrolled in the study . Mean age, gender distribution, and chronic pain duration were comparable between the groups . Unexpec2 (147) = 4.034, p = 0.058]. The results showed that overall pain intensity decreases after rTMS stimulation as indicated by the significant main effect of factor time  = 194.14, p ≤ 0.001, d = 0.961) and interaction of time and session  = 3.73, p = 0.027, d = 0.440)  = 4.77, p = 0.010, d = 0.527). Overall pain intensity on the 4th week  of stimulation was comparable to baseline (57.72%), while pain intensity on the 36th week of stimulation was significantly lower  than baseline. The main effect of the factor group was only nearly significant  = 3.99, p = 0.009, d = 0.457). Bonferroni corrected post hoc t-test showed that at baseline and 4th week after stimulation, there were no significant differences in overall pain intensity between the groups (p ≤ 0.001) than in the M1-group (56.11%) (t-test showed that the significant differences between the DLPFC- and M1-group’s pain intensity on the 36th week was driven by the significantly lower (p ≤ 0.001) pain intensity in the former (47.83%) than the latter (75.81%) before stimulation (p = 0.085).For the GPQ-NPRS, we interpreted a full model because the Pearson Chi-Square test indicated the goodness of fit of this model to the data , anxiety score , and stress score . For the depression score, the model revealed no significant results  = 6.90, p = 0.003, d = 0.433). Bonferroni corrected post hoc t-test showed a significant decrease (p = 0.001) in anxiety score on the 4th week (4.09) of stimulation compared to baseline (6.47). This indicates a change of overall anxiety level from moderate to mild  = 12.19, p ≤ 0.001, d = 0.315). Bonferroni corrected post hoc t-test showed a significant decrease in stress score on the 4th week  and 36th week  compared to baseline (10.17). These results indicate a change in overall stress level from moderate at baseline to normal on the 4th and 36th week of stimulation (Pearson Chi-Square test results indicated goodness of fit of the models for the depression score  and MCS  data were violated. Thus, we modelled the two data sets together. In this full model, the between-subject factor “group”  and the within-subject factor “session” , and “composite summary”  are the fixed factors. The goodness-of-fit of the full model is better than individual models . The results of the full model showed a significant main effect of group  = 3.75, p = 0.026, d = 0.213). Bonferroni corrected post hoc t-test indicated a significantly higher (p = 0.016) overall score in the DLPFC-group (40.47) than the sham group (35.06). The main effect of session was also significant  = 3.36, p = 0.002, d = 0.456) as indicated by the significantly higher overall score in the 4th  and 36th week  compared to baseline (35.87). Similarly, the main effect of composite summary was significant  = 61.65, p ≤ 0.001, d = 0.642) as indicated by the significantly higher (p ≤ 0.001) MCS (44.89) than PCS (31.86) (t-test showed that the significantly higher (p = 0.001) MCS  = 2.46, p = 0.089, d = 0.241).For the SF-12 questionnaire, we initially modelled PCS and MCS, separately. However, Pearson Chi-Square test indicated that the goodness of fit of this model to the PCS [X (31.86) A,B. Addi001) MCS B in the This study investigated the potential of multi-session high-frequency rTMS over M1 and DLPFC to treat chronic pain patients. To our knowledge, this is the first investigation of analgesic effects of rTMS applied to these brain regions using an MRI-guided neuronavigation method in a single study. In the active treatment groups (M1 and DLPFC-group), we monitored changes in the sensory-discriminative aspect of pain utilising the GPQ and affective/emotional aspect using the DASS and SF-12 questionnaires for 36 weeks. Assessment in the Sham-group ended after 4 weeks. Compared to left M1 stimulation, the results showed that 36 weeks of left DLPFC stimulation could reduce pain perception and improve health-related quality of life.In the earlier periods of stimulation, pain perception is reduced as indicated by the reduction in pain intensity scale at baseline and on the 4th week post-stimulation, as well as by the decrease in the AUC covering the first five days of stimulation. However, the reduction in pain perception was comparable between the three groups, which can be due to a robust placebo effect in the sham group or minimal effect of active stimulation in the M1 and DLPFC-group. Robust placebo effect cannot be rule out since a non-significant trend in pain reduction is evident in the sham group in the early period of the study. Placebo effect is common in pain therapies and is thought to arise from expectancy-induced analgesia ,48. For The possibility of M1 and DLPFC stimulation having minimal effects on pain perception during the early sessions, on the other hand, find support from a meta-analysis of pain studies showing that high-frequency multiple-dose rTMS  only had minor short-term effects on chronic pain . The autn = 28) constituted a relatively larger sample size than in our group (n = 11). Alternatively, the effectiveness of rTMS for pain relief can be influenced by pain chronicity [Concerning the effect of stimulation in later sessions, we observed that pain perception reduction is lower in magnitude in the M1-group than the DLPFC-group. For the M1-group, the weaker effect was persistent since pre- and post-stimulation pain perception at baseline, 4th and 36th week has a waxing and waning pattern A, indicaronicity . Patientronicity .For the DLPFC-group, pain perception reduction was robust compared to that of the M1-group at later sessions. This was indicated by the significantly lower pain intensity scale compared to baseline and compared to those of the M1 group at the same time point. In the DLPFC-group, although post-stimulation pain perception at baseline, 4th and 36th weeks were comparable, there was an evident decline in the pre-stimulation pain perception suggesting that pain no longer reverts to baseline level after each session B. There Concerning the modulation of pain’s affective/emotional aspect by rTMS, the DASS survey only showed overall improvement in anxiety (moderate to mild) and stress . A significant change in anxiety level is observed between baseline and the 4th week without group-specificity. For stress, the significant decrease on the 36th week compared to baseline is only driven by the M1 and DLPFC-group since there were no measurements for the sham-group at this period. Although the differences between the groups did not differ statistically, it was evident that the mean stress level on the 36th week was lower in the DLPFC group (6.50) than the M1-group (8.50), indicating normal and mild stress levels, respectively. Improvement in stress level was only present in the DLPFC group, from moderate at baseline to normal on the 36th week. In contrast, in the M1-group, stress levels did not change from baseline to 36th week (both mild). Modulation of brain structures linked to the affective/emotional aspect of pain, such as the cingulate cortex through cortico-subcortical pathways, can directly account for stress level improvement in the DLPFC-group. Imaging studies provided evidence that left DLPFC rTMS also affects blood flow and metabolism in the ACC . The ACCThe results of the present study indicate that multi-session rTMS of the left DLPFC leads to significant improvement in pain perception and stress level reduction. These effects are better than those obtained from left M1 stimulation, where no effective pain relief was elicited. This indicates an advantage of the DLPFC as a target area for pain rehabilitation by multi-session rTMS. However, the following limitations of our study must be taken into account. First, there were no measurements from the sham group at later sessions. We considered this a significant drawback of the study because comparisons in those time points are only limited between the M1- and DLPFC-group. In our opinion, data comparisons are not entirely non-trivial because the sham and M1 stimulation (both stimulated left M1) have comparable effects at early time points, while the comparison of data from two separate brain areas (M1 vs. DLPFC) finally revealed significant differences. Second, our sample size was relatively small; hence, further studies with a larger population are warranted. Finally, the patients’ maintenance medications  and analgesics) were not discontinued during the study. There are reports that analgesics  affects cortical excitability . At the


These findings pave the way for the scalable synthesis of crystalline graphdiyne-based materials as cohesive thin films.Graphdiyne polymers have interesting electronic properties due to their π-conjugated structure and modular composition. Most of the known synthetic pathways for graphdiyne polymers yield amorphous solids because the irreversible formation of carbon–carbon bonds proceeds under kinetic control and because of defects introduced by the inherent chemical lability of terminal alkyne bonds in the monomers. Here, we present a one-pot surface-assisted deprotection/polymerisation protocol for the synthesis of crystalline graphdiynes over a copper surface starting with stable trimethylsilylated alkyne monomers. In comparison to conventional polymerisation protocols, our method yields large-area crystalline thin graphdiyne films and, at the same time, minimises detrimental effects on the monomers like oxidation or cyclotrimerisation side reactions typically associated with terminal alkynes. A detailed study of the reaction mechanism reveals that the deprotection and polymerisation of the monomer is promoted by Cu( We present a one-pot deprotection/polymerisation protocol for the synthesis of crystalline graphdiynes on top of a copper surface starting with stable trimethylsilylated alkyne monomers.  In recent years, a triazine (C3N3)-based, two-dimensional π-conjugated carbon nitride (TGCN) with a narrow band-gap was identified as a “beyond graphene” material.2,3 However, to this day TGCN remains an exotic and singular addition to the graphene family because its synthetic protocol requires high-temperature, high-pressure, and ionothermal conditions, and this type of linking chemistry does not readily translate to other, suitable molecular building blocks. Graphdiynes are a class of materials that satisfy two requirements for “post graphene” materials: (i) structurally, graphdiynes are two-dimensional, layered, and periodic,4 (ii) chemically, they are covalently linked, fully π-conjugated, and modular in their make-up.5 In a periodic graphdiyne polymer, all carbon atoms are sp2–sp hybridised. This enables charge-transport across the π-conjugated framework and is important in the development of future carbon-based electronic devices.6–10 Despite promising conceptual attributes, the synthesis of highly ordered, defect-free graphdiynes remains a challenge as most synthetic protocols yield amorphous powders.11–15 To address this, Glaser-type coupling, Eglinton coupling, and Glaser–Hay coupling have been proposed, where the oxidative homocoupling of terminal alkynes is achieved in the presence of dissolved copper salts or copper surface under alkaline conditions.11,16,17 However, the free rotation around aryl–alkyne single bond and detrimental oxidation and self-polymerization side reactions of the terminal alkynes have so far resulted in the formation of less ordered graphdiynes polymers.6,9,18–20 Additionally, the kinetics of the underlying coupling reactions determine the morphology of the graphdiyne polymer. We found that faster Sonogashira coupling yields amorphous graphdiyne polymers, while the more sluggish Glaser protocol leads to ordered, crystalline polymer frameworks.21 Alternatively, a modified Hiyama protocol has recently been introduced for the synthesis of graphdiyne polymers. Here, trimethylsilyl-protected alkynes are deprotected in situ and polymerise to graphdiynes. However such protocols use dissolved copper salts and or multi-component catalytic systems which adds to the chemical complexity which has so far limited the mechanistic understandings.22,23 Previously, we showed that monomers with terminal alkyne functional groups can be polymerised at the liquid–solid interface using a copper foil as a physical template and promoter of Glaser–Hay coupling and that this reaction yields continuous polymer films that grow on the copper surface. However, we found that during the initial stages of this reaction, the relatively high concentration of alkyne monomers at the copper interface favours a [2 + 2 + 2] cyclotrimerisation of surface-bound species.6 To address the shortcomings of conventional synthetic protocols for graphdiyne polymers, we propose to combine the Glaser and the modified Hiyama coupling. In the following, we obtain crystalline triazine-based graphdiyne (TzG) films directly from 2,4,6-tris(4-[(trimethylsilyl)ethynyl]phenyl)-1,3,5-triazine (TMS-Tz) monomers over a commercial copper foil surface. Where the copper surface (i) facilitates slow deprotection and fast coupling of the trimethylsilylated alkynes, (ii) acts as a template for the polymer growth which favors in-plane coupling of alkynes, and (iii) prevents the cyclotrimerization side-reactions typically associated with terminal alkynes. The resulting graphdiyne films have a thickness of up to 3 μm and lateral dimensions of 2–3 cm. In this study, we also shed light on the mechanism of monomer coupling. We observe that the copper oxide/hydroxide layer on as-received copper foil promotes simultaneous deprotection and polymerisation of trimethylsilylated alkynes  was dissolved in pyridine, and a commercial copper foil was immersed in the solution as a physical template and initiator of the coupling reaction. The reaction mixture was then stirred at 60 °C for 72 h, which afforded 1,4-diphenylbuta-1,3-diyne (Ph–CC–CC–Ph) as the only product in quantitative yields (via slow deprotection of the trimethylsilyl (TMS) groups and subsequent coupling to yield the diyne bridge, which was confirmed by gas chromatography  molecule.Inspired by the recently introduced, modified Hiyama coupling and our previous know-how in surface-assisted synthesis, we investigated the polymerisation of alkynyl silanes to graphdiyne polymers. In a control experiment, trimethylsilylethynylbenzene  on a copper surface directly using 2,4,6-tris(4-[(trimethylsilyl)ethynyl] phenyl)-1,3,5-triazine (TMS-Tz) monomers dissolved in pyridine, to gain control over the structure and morphology we optimized the reaction condition by varying the reaction temperature , breathing (at 1359 cm−1), and out-of-plane ring bending (at 816 cm−1) modes of the triazine ring are retained.21,25,26The chemical composition of isolated TzG flakes was confirmed by combustion elemental analysis after drying at 180 °C for 72 h as 83.50 wt%, 4.04 wt% (H), and 10.44 wt% (N), compared to theoretical values of 85.73 wt% (C), 3.17 wt% (H), and 11.11 wt% (N) Table S1. Thermogy of TzG  and S16†C–CC–) bridges was found at 2200 cm−1 by Raman spectroscopy  magic-angle spinning (MAS) solid-state nuclear magnetic resonance (NMR). We observe the presence of diyne bridges at 82 and 77 ppm. The triazine ring carbon environment is assigned to the peak at 170 ppm, and the phenyl carbon environments are assigned between 128–140 ppm ; these reflections correspond to the (100), (200), and (220) diffraction planes.Powder X-ray diffraction data and electron diffraction pattern show the formation of a crystalline TzG framework . We mode6,21 Hence, there is a high likelihood of stacking defects and turbostratic disorder between neighbouring layers. As a consequence, pore channels are occluded and the sample appears to have no accessible surface area (by N2 BET at 77 K).According to DFT calculations there is only a very low energy barrier (of up to 0.5 eV per unit cell) between the different stacking modes of TzG.28,29 We estimated the thickness of TzG films from SEM images taken parallel to the basal plane of the layers as approx. 3 μm. The low scattering contrast and small irradiated sample volume of TzG films give rise to fairly poor information content in PXRD profiles. Thus, we performed high-resolution transmission electron microscopy (HR-TEM) and selected area electron diffraction (SAED) measurements to elucidate the microstructure of TzG films  was used to study the morphology and layer thickness of the TzG  and S20†The optical properties of the synthesized TzG films were then evaluated by solid-state UV/vis diffuse reflectance spectroscopy and solid-state photoluminescence (PL) spectroscopy. UV/vis spectra show that TzG has a discernible absorption edge at 480 nm Fig. S23a, correspii) species that are reductively eliminated as Cu(i) during the formation of the diyne-bridge. The re-oxidation of Cu(i) to Cu(ii) is then achieved by oxygen  species found on the surface of the as-received copper foil. Therefore, to better understand the role of surface species in the proposed deprotection/polymerisation cascade, we performed series of control experiments.Conventional Glaser coupling of alkynes uses stoichiometric amounts of a copper we attempted to grow TzG using CuO under ambient atmospheric conditions  acetylide complex is formed. This complex undergoes reductive elimination of Cu(i) during the formation of the butadiyne bridge. In this mechanism, trimethylsilanols (Me3Si–OH) observed via GCMS act as a proton source for the deprotected alkynes (In control experiment (1) we prepared a clean Cu(0) surface by argon ion sputtering. XPS analysis confirmed the removal of any oxides present on the surface , S26†. N alkynes .ii) satellite signals between 940–945 eV and the main Cu 2p3/2 signal at 934.9 eV, unambiguously confirming the presence of the native Cu(ii) oxides.31 After TzG growth, the main Cu 2p3/2 component shifts to 932.8 eV which can either be assigned to Cu(0) or Cu(i) oxidation states.31,32 Thus, to differentiate between these two species, we compared the main Cu 2p3/2 component of the TzG covered Cu foil with the argon-ion sputtered copper foil, which showed a peak at a slightly lower energy of 932.6 eV. Since we encountered charging of TzG during XPS measurements, it is difficult to solely rely on the binding energies to determine the oxidation state of the probed element. Therefore, to elucidate the oxidation state of the copper foil after polymerisation, we measured the Cu LMM Auger region of the same copper surfaces (i) in the as-received state, (ii) after TzG growth, and (iii) after argon ion sputtering. The Auger electron spectra of the TzG coated copper foil show a broad and asymmetrical peak centered at 916.1 eV, indicating the presence of mainly Cu(i) species . These findings substantiate the proposed mechanism  in as-received state, (ii) after TzG growth, and (iii) after argon ion sputtering . The Cu  species  and S30†echanism .In summary, we present a simple, one-pot protocol for the synthesis of crystalline triazine-based graphdiyne films with a thickness of up to 3 μm and lateral dimensions of 2–3 cm directly from trimethylsilylated alkynes on commercial copper foil. In this work, we have merged the concepts of on-surface Glaser coupling with the ease-of-use of Hiyama coupling into a deprotection/polymerisation reaction. In principle, the here-presented method can be easily extended to a plethora of suitable monomer building blocks for the synthesis of freestanding, ordered, diyne-bridge polymer films.The datasets supporting this article have been uploaded as part of the ESI.†All authors have actively contributed to the work presented in this paper R. K. and M. J. B. conceived the research project. R. K. planned and carried out the synthetic experiments, analysed the data and wrote the paper. J. H. carried out the XPS measurements and analysis. M. T. carried out the mechanistic studies and assisted in manuscript writing. D. B. carried out the argon plasma experiments and analysis. J. M. carried out the SEM and SEM-EDX measurements. P. A. performed the XPS measurements and Wagner plots. A. M. helped with the glove box experiments. M. D. performed the solid-state NMR experiments. D. K optimized the GC method and did the GCMS measurements and M. J. B. conceived the experiments helped in correcting the paper.There are no conflicts to declare.SC-012-D1SC03390E-s001